Abstract

Background Longitudinal studies have been in conclusive in
identifying alcohol as a risk factor for anxiety and depression.

Aims To examine whether excessive alcohol consumption is a risk
factor for anxiety and depression in the general population, and whether
anxiety and depression are risk factors for excessive alcohol consumption.

Method Data were analysed from the 18-month follow-up of the
Psychiatric Morbidity Among Adults Living in Private Households, 2000
survey.

Results Hazardous and dependent drinking were not associated with
onset of anxiety and depression at follow-up. Binge-drinking was
non-significantly associated with incident anxiety and depression (adjusted
OR=1.36, 95% CI 0.74-2.50). Abstainers were less likely to have new-onset
anxiety and depression at follow-up. Anxiety and depression or sub-threshold
symptoms at baseline were not associated with incident hazardous or
binge-drinking at follow-up, but there was weak evidence linking sub-threshold
symptoms with onset of alcohol dependence (adjusted OR=2.04, 95% CI
0.84-4.97).

Conclusions Excessive alcohol consumption was not associated with
the onset of anxiety and depression but abstinence was associated with a lower
risk. Sub-threshold symptoms were weakly associated with new-onset alcohol
dependence.

The Alcohol Harm Reduction Strategy for England
(Prime Minister’s Strategy Unit,
2004) outlines a range of measures to reduce the public’s
consumption of alcohol. Heavy alcohol consumption has been implicated in the
development of anxiety and depression
(Schuckit, 1983). Many
cross-sectional studies have identified considerable comorbidity between
anxiety and depression, and alcohol abuse. For example, data from four large
community-based epidemiological studies (nÏ 22 000) in Europe
and the USA consistently demonstrated a two- to threefold increase in the
lifetime prevalence of anxiety and depression in those with DSM–III or
DSM–III–R alcohol abuse or dependence
(Swendsen et al,
1998). The temporal nature of the association is difficult to
determine from cross-sectional studies, with uncertainty arising as to whether
alcohol is a risk factor or a form of self-medication
(Miller et al, 1996).
The results of existing longitudinal studies on the relationship between
alcohol consumption and anxiety and depression are conflicting. An early
meta-analysis of eight longitudinal studies found that baseline alcohol
consumption was significantly associated with later depression
(Hartka et al, 1991).
However, little adjustment for confounding was made. More recent reports have,
in general, found no association between alcohol consumption and incident
depressive illness (Moscato et al,
1997; Wang & Patten,
2001), although there is some evidence that women may be at
greater risk (Wang & Patten,
2001). Gilman & Abraham
(2001) observed increased odds
of major depression at 1 year when diagnosed with alcohol dependence at
baseline, also finding that women were at greater risk. The 18-month follow-up
of participants of the Psychiatric Morbidity Among Adults Living in Private
Households, 2000 survey (Singleton &
Lewis, 2003) provides an opportunity to determine whether
excessive alcohol consumption and abnormal patterns of use are risk factors
for incident anxiety and depression in the general population. To our
knowledge this is the first such study performed in England and Wales. The
study also examined the reverse relationship, considering whether anxiety and
depression are risk factors for the development of abnormal patterns of
alcohol consumption.

METHOD

Psychiatric Morbidity Among Adults Living in Private Households,
2000

Data were used from the 18-month followup of the Psychiatric Morbidity
Among Adults Living in Private Households, 2000 survey
(Singleton & Lewis, 2003).
The original study was a cross-sectional survey of a nationally representative
sample of 8580 adults (aged 16–74 years) living in private households in
Great Britain (Singleton et al,
2001). Participants in the original survey were classified
according to their score on the Clinical Interview Schedule – Revised
(CIS–R; Lewis et al,
1992). All those identified as having a mental disorder
(CIS–R score ≥ 12, probable psychosis, or drug or alcohol dependence)
at the time of the cross-sectional survey and those with sub-threshold
symptoms (CIS–R score 6–11) were eligible for follow-up. In
addition, a random 20% of those with no evidence of a mental disorder
(CIS–R score Î 6) were also followed-up.

In total, 3536 participants were selected for follow-up. Of these, 3045
were located by the interviewers. For the remaining 491 (14% of the sample)
contact was not possible because the household had moved and could not be
traced, or for other reasons, such as the death of the individual. Of the 3045
who were successfully located, 2413 (79%) completed the follow-up interview,
503 (17%) refused to be interviewed, and for 129 (4%) the interviewer was
unable to contact the person.

Ethical approval for the study was obtained from the London Multi-Centre
Research Ethics Committees in England.

Measurement of alcohol use

Most information was collected face-to-face by lay interviewers using
computer-assisted interviewing. However, responses for questions about alcohol
and drug use were directly entered into the computer by the participants
themselves.

In both the baseline and the 18-month follow-up surveys, alcohol use was
recorded using the Alcohol Use Disorders Identification Test (AUDIT;
Saunders et al, 1993).
The AUDIT comprises ten questions relating to alcohol use and its consequences
in the previous 12 months. A score of 8 or more out of 40 has been suggested
to denote hazardous alcohol use (Saunders
et al, 1993).

Those who scored 10 or more on the AUDIT were asked to complete the
Severity of Alcohol Dependence Questionnaire (SAD–Q;
Stockwell et al,
1983) to assess dependence. The SAD–Q consists of 20
questions, covering a range of symptoms of dependence, each scored from 0 to
3. The reference period is the 6 months prior to the interview. A total score
of 3 or less indicates no dependence, a score of 4–19 indicates mild
dependence, 20–34 indicates moderate dependence and 35–60
indicates severe dependence.

Alcohol use was classified in four ways:

hazardous drinking: AUDIT score ≥ 8;

above government guidelines: more than 21 units per week for men or more
than 14 units per week for women;

binge-drinking: six or more drinks on one occasion on at least a monthly
basis (same definition used for men and women);

dependence: AUDIT score ≥ 10 and SAD–Q ≥ 4.

Alcohol use above government guidelines was based on two AUDIT questions
concerning the frequency and amount of alcohol consumed
(Table 1). Those classified as
exceeding guidelines were identified using the following combinations of
responses: men – A4 B5, A5 B4, A5 B5 and women – A4 B4, A4 B5, A5
B3, A5 B4, A5 B5.

Ascertainment of alcohol use above government guidelines using the
Alcohol Use Disorders Identification Test for frequency and quantity

At baseline the AUDIT assessed use over the year prior to interview; at the
follow-up interview the reference time period related to the whole period
between interviews. The SAD–Q assessed dependence in the 6 months prior
to interview in both surveys.

Measurement of psychiatric morbidity

Anxiety and depression was used as a diagnostic category, as most people
with significant psychiatric problems have symptoms of both, and many meet the
criteria for more than one diagnosis. The CIS–R has been validated as a
measure of common mental disorders (Lewis
et al, 1992), covering diagnoses of depressive illness,
generalised anxiety disorder, obsessive–compulsive disorder, panic
disorder, phobias, and mixed anxiety and depressive disorder. It comprises 14
sections, with possible scores within each ranging from 0 to 4 (except the
section on depressive ideas which has a maximum score of 5). A total score of
12 or more was used to indicate the presence of disorder. Owing to questions
relating only to the previous week, a true measure of incident anxiety and
depression was not obtainable for the period between baseline and follow-up,
as cases may have presented and then subsequently recovered. However, this
phrase, or ‘new onset’, will be used as shorthand with the
understanding that a random misclassification may have occurred. It is
recognised that the CIS–R can be used to diagnose generalised anxiety
disorder, and yet produce a score less than 12. This occurred in very few
cases at baseline, and so the ensuing degree of bias was small.

Data-set

In total, 2406 participants completed the baseline and follow-up surveys.
Of these, 750 had a CIS–R of 12 or more at baseline and were therefore
excluded from analyses examining predictors of anxiety and depression at 18
months’ follow-up. The cohort therefore comprised 1656 individuals, of
whom 1578 (95%) had data available on a range of potential confounders at
baseline (indicators of socio-economic status, life events, type of area
(urban/rural), size of primary support group, current smoking habits, illicit
drug use in the previous year, use of psychotropic drugs or therapy, hospital
treatment in the past 3 months for mental health problems, and consultations
with mental health professional(s) in the past year).

Statistical analysis

All analyses were conducted using Stata version 8
(Stata Corporation, 2003).
Probability weights were used to account for the stratified sampling procedure
and non-response in all analyses.

Logistic regression was used to examine the association between alcohol use
and onset of anxiety and depression (CIS–R score 512) at 18 months.
Univariate associations (in terms of odds ratios and their 95% CI) are
reported. Associations were adjusted for baseline CIS–R score and
potential confounding factors, both individually and cumulatively.

Further analyses examined the association between anxiety and depression at
baseline and alcohol use (binge-drinking, hazardous drinking or dependence) at
follow-up. Individuals who were classified as binge drinkers (nd
752), hazardous drinkers (nd 669), or dependent on alcohol
(nd 309) at baseline were excluded from these additional analyses.
Complete data, including information on possible confounders, were available
for 1562, 1645 and 1987 individuals, respectively.

RESULTS

After weighting to account for the stratified sampling strategy and
non-response, the prevalence of hazardous drinking was 24% at baseline
(Table 2). Only 6% of the
population reported drinking in excess of government guidelines, but the
prevalence of binge-drinking was substantially higher (31%).

Eighteen per cent of the population reported binge-drinking at least once
per week; 7% of the population were dependent on alcohol
(Table 2); 11% of the
population reported abstinence from alcohol over the preceding 12 months
(Table 2). Overall, alcohol use
was more prevalent among men (Table
2); 41% of men reported monthly binges compared with 21% of women.
Men were almost six times more likely to be dependent on alcohol (weighted
prevalence 11.6%) than women (2.0%; Table
2).

Associations between baseline alcohol consumption and anxiety and
depression at follow-up

Of the 1656 individuals who were not classified as having anxiety and
depression at baseline, 184 had a CIS–R score of 12 or more at follow-up
(weighted prevalence 6.3%, 95% CI 5.0–7.6). Hazardous drinkers (AUDIT
score ≥ 8) did not have an increased odds ratio of developing anxiety and
depression at follow-up compared with non-hazardous drinkers (adjusted odds
ratio=0.76, 95% CI 0.42–1.36). Those who had not consumed alcohol in the
preceding 12 months were less likely to develop anxiety and depression at
follow-up compared with non-hazardous drinkers. This association strengthened
after adjustment for baseline CIS–R and potential confounders (adjusted
odds ratio=0.36, 95% CI 0.17–0.77). Those individuals who drank above
government guidelines at baseline had a comparable odds of anxiety and
depression at followup as those who drank within recommended limits (adjusted
odds ratio=0.87, 95% CI 0.43–1.74)
(Table 3).

Associations between alcohol consumption and anxiety and depression at
follow-up

Unadjusted analyses suggested that those who reported binge-drinking at
least once per month were more likely to develop anxiety and depression at
follow-up than non-binge drinkers (odds ratio=1.58, 95% CI 0.97–2.56).
However, when adjusted for baseline CIS–R and potential confounders,
this association was attenuated (odds ratio=1.36, 95% CI 0.74–2.50).
Stratifying drinkers according to the frequency of binge-drinking provided
little evidence for a dose–response relationship
(Table 3).

Those classified as dependent on alcohol at baseline (AUDIT ≥10 and
SAD–Q ≥4) had an increased likelihood of anxiety and depression at
follow-up (unadjusted odds ratio=1.61, 95% CI 0.91–2.87), although this
association was not statistically significant. Again, this association
attenuated when adjusted for baseline CIS–R and other confounders
(adjusted odds ratio=1.09, 95% CI 0.55–2.17).

Irrespective of the method used to classify alcohol consumption, those who
had not drunk alcohol in the previous 12 months were significantly less likely
to have anxiety and depression at follow-up
(Table 3).

Stratifying by gender showed some differences
(Table 4). Men who binged at
least once per month had a threefold increased risk of anxiety and depression
at follow-up after adjustment for confounders. In contrast, no excess was
observed for female binge drinkers. However, a test for interaction did not
provide statistical support (P=0.30).

Gender-specific associations: alcohol consumption and anxiety and
depression at follow-up

Associations between baseline anxiety and depression and alcohol
consumption at follow-up

There was no excess of monthly binge-drinking at follow-up in those with
sub-threshold symptoms (CIS–R score 6–11) or anxiety and
depression (CIS–R score ≥12) at baseline (adjusted odds ratio=1.04,
95% CI 0.58–1.84 and 0.95, 0.51–1.80, respectively;
Table 5), nor was there an
excess of hazardous drinking in these groups (adjusted odds ratio=1.27, 95% CI
0.76–2.12 and 1.05, 0.53–2.07, respectively). However, those with
CIS–R scores above 5 were almost twice as likely to develop alcohol
dependence at followup as those with lower scores (CIS–R score
0–5), although this failed to reach statistical significance
(Table 5).

Associations between anxiety and depression and alcohol use at
follow-up

Again, stratification by gender showed some differences
(Table 6). Men with
sub-threshold symptoms or anxiety and depression at baseline had approximately
a twofold increased odds of binge-drinking at follow-up. In contrast, women
with anxiety and depression had a reduced odds of binge-drinking at follow-up.
However, this interaction was not statistically significant (P=0.23).
There was evidence that men with anxiety and depression and women with
sub-threshold symptoms at baseline had an increased odds of alcohol dependence
at follow-up. This bordered on statistical significance (P=0.07).

Gender-specific associations: anxiety and depression and alcohol
consumption at follow-up

DISCUSSION

This study aimed to determine whether excessive alcohol consumption, and
abnormal patterns of alcohol use were risk factors for ‘incident’
anxiety and depression. Data were analysed from the longitudinal follow-up of
participants in the Psychiatric Morbidity Among Adults Living in Private
Households, 2000 survey (Singleton &
Lewis, 2003).

Findings

Hazardous drinking, as defined by an AUDIT score of 8 or greater, was not
associated with incident anxiety and depression at follow-up. Binge-drinking
(on at least a monthly basis) was associated with an excess of anxiety and
depression, but this did not reach statistical significance. After adjustment
for confounders, there was no association between dependent drinking (AUDIT
score ≥10 and SAD–Q ≥4) and onset of anxiety and depression at
followup. Those who had not consumed alcohol within the previous 12 months
consistently had a reduced odds of developing anxiety and depression.

Analyses stratified by gender suggested that men who binge drank (on at
least a monthly basis) had a threefold increased odds of anxiety and
depression at followup compared with men who did not binge drink. No such
association was observed for women.

The reverse analysis did not demonstrate an excess of hazardous or
binge-drinking at follow-up in those with anxiety and depression at baseline.
There was some evidence that men with sub-threshold symptoms or anxiety and
depression at baseline had an increased odds of binge-drinking at follow-up,
although this gender differential was not statistically significant. Those
with sub-threshold symptoms or anxiety and depression at baseline had a
twofold increased odds of reporting alcohol dependence at follow-up. This was
of borderline significance. Stratification by gender demonstrated that men
with anxiety and depression at baseline had a twofold increased odds of
alcohol dependence at follow-up, whereas women with sub-threshold symptoms had
a fivefold increased odds of dependence. The test of interaction bordered on
statistical significance.

Comparison with previous longitudinal studies

Our findings are in direct contrast to an early meta-analysis
(Hartka et al, 1991)
that reported a significant correlation between baseline consumption of
alcohol and depression at follow-up based on data from eight longitudinal
studies. However, in this analysis control of confounders was limited to age,
gender and interval between measurements.

Our findings are also in direct contrast to those of Gilman & Abraham
(2001) who observed that after
adjustment for confounders, including baseline depression score, age,
socio-economic status, ethnicity, and site, women (odds ratio=3.52) and men
(odds ratio=1.77) who were dependent on alcohol at baseline had a
significantly increased odds of major depression according to DSM–IV
criteria (American Psychiatric Association,
1994) after 1 year of followup. The odds ratio for men falls
within the 95% CI calculated for the present study, although that for women
does not.

Overall, our findings were consistent with those of Wang & Patten
(2001) who analysed a
longitudinal cohort of the Canadian National Population Health Survey,
numbering 11 000. Taking DSM–IV major depression as an end-point, rather
than the criteria used in our study, they observed no excess morbidity among
those who drank daily, those who drank in binges (more than five drinks),
those who had more than one drink daily, and among drinkers in general.
Alcohol dependence was not considered. Similarly, in a randomly selected
community cohort with follow-up at 3 and 7 years, Moscato et al
(1997) found no excess
incidence of depressive symptoms among those with ‘alcohol
problems’ (defined as a DSM–IV diagnosis of alcohol dependence or
abuse or drinking more than five drinks a day on one or more occasions per
week).

Wang & Patten (2001)
reported an excess of major depression in binge-drinking women compared with
non-binge drinkers. This contrasts with our finding that men who binge drank
had an increased odds of anxiety and depression at followup. These
gender-specific differences are difficult to interpret given the differences
in definitions of alcohol consumption and psychiatric morbidity.

Moscato et al
(1997) also performed the
reverse comparison and noted that depressive symptoms were associated with
incident alcohol ‘problems’ in women but not in men. After
adjustment for confounders, this effect among women was noted to be stronger
in the short (3 years) rather than longer term (7 years). In our study,
subclinical anxiety and depression (CIS–R 6–11) was associated
with an increased odds of alcohol dependence at followup, with evidence that
this effect was stronger among women, although confidence intervals were wide.
For this reason, interpretation of the gender-specific estimates must be
viewed with caution.

Our findings are partially consistent with those of Lipton
(1994). Data from the Los
Angeles Epidemiological Catchment Area study showed evidence for a U-shaped
relationship between alcohol use at time 1 and depressive symptomatology 1
year later, in the presence of financial strain and negative life events;
heavy drinking, when compared with light–moderate or moderate drinking,
was associated with a 50% increase in depressive symptom score. However, a
similar increase in depressive symptom score was observed when abstinent or
light drinking was compared with light–moderate or moderate drinking.
This effect was attenuated in the absence of negative events or financial
strain. This association was only observed in the sub-group analysis; when all
participants were included in the analysis, a suggestion of a U-shaped
relationship remained, but with confidence intervals that did not support a
difference between drinking groups. Their finding of increased risk among
non-drinkers is at odds with the consistent finding in the present study of
reduced risk among non-drinkers. The difference in findings may be because the
present study considered more possible confounding variables, including those
relating to previous mental illness and ongoing psychotropic therapy. Without
consideration of such variables the estimate of the association might have
been biased by non-drinkers who were abstinent owing to previous mental
illness and were at risk of relapse, therefore giving a false impression of
the risk/benefits associated with abstinence.

Strengths and limitations

By using national population data, we have avoided the selection and
referral biases inherent in studies of clinic-based patients. Observer bias
was eliminated by the participants themselves entering data directly onto a
laptop computer. The study design reduced the chance of recall bias, as might
be found in a retrospective case–control study. In addition, well
validated tools were used to measure alcohol consumption (AUDIT and
SAD–Q; Stockwell et al,
1983; Saunders et al,
1993) and anxiety and depression (CIS–R;
Lewis et al,
1992).

The definition of drinking above government guidelines was calculated using
two items from the AUDIT. A direct measure of the quantity of alcohol consumed
was lacking and it is probable that our figures are an underestimate. Indeed,
prior community surveys have estimated that 27% of men and 15% of women drink
above government guidelines
(http://www.performance.doh.gov.uk/hpsss/tbl_a9.htm),
which is substantially higher than our estimates (9 and 3%, respectively).
Asking participants to recall the amount of alcohol consumed over a shorter
period (for example 1 week) would provide a more accurate estimate of the
alcohol consumed, although the representativeness of such data may be
questioned.

Although a comparatively large number of individuals were surveyed in this
study, the power to detect associations for alcohol dependence, in particular,
was limited. The possibility of a type II error remains. Others have also
commented on their limited ability to detect associations for particular
patterns of drinking (Wang & Paten,
2001).

There is a high probability of random misclassification, owing to the true
incidence over the follow-up period not being obtained, but rather a snap-shot
picture of mental health for the week prior to follow-up. It is therefore
possible that cases of anxiety and depression might have emerged and
subsequently recovered, and therefore might not be counted. This random
misclassification would affect all participants, but would make a
statistically significant result less likely.

This study (as others; Wang &
Patten, 2001) has a relatively short follow-up period (18 months).
A longer period of follow-up might have resulted in significantly more at-risk
drinkers developing anxiety and depression. This is particularly relevant, as
the analysis adjusted for baseline CIS–R and AUDIT scores. The
association between baseline alcohol consumption and onset of anxiety and
depression, and vice versa, might have been underestimated by correcting for
baseline CIS–R and AUDIT scores, which as measures of subclinical
disease may be on the causal pathway. However, unadjusted data showed little
evidence of a significant association, and therefore it is unlikely that
overadjustment is the sole explanation for the lack of association.

Anxiety and depression and alcohol consumption

In summary, hazardous drinkers did not have increased odds of anxiety and
depression at follow-up; there was a suggestion that binge-drinking and
dependence are risk factors for anxiety and depression, but sample size was
insufficient for firm conclusions. However, those who abstained from alcohol
had a reduced risk. Participants with sub-threshold symptoms or anxiety and
depression at baseline had increased odds of reporting alcohol dependence at
18 months; this bordered on statistical significance.

Public health implications

The protective effect of abstinence compared with an ‘
acceptable’ drinking pattern is most notable. This suggests that
a ‘safe’ level of drinking (in terms of the prevention of anxiety
and depression) may be lower than previously recognised. The recent
publication of the Alcohol Harm Reduction Strategy for England
(Prime Minister’s Strategy Unit,
2004) sets out more conservative guidelines for drinking:
3–4 units per day for men and 2–3 units per day for women, but
such guidelines were devised to reduce both the social and physical problems
associated with excessive alcohol consumption, not only the risk of anxiety
and depression. Further work is therefore required to examine the effect of
drinking at a lower threshold on the risk of anxiety and depression before
guidance can be provided.

Clinical Implications and Limitations

CLINICAL IMPLICATIONS

Abstinence protects against the development of anxiety and depression.

Binge-drinking might be a modifiable risk factor for anxiety and
depression, especially among men.

Subclinical anxiety and depression is suggested as a risk factor for the
development of alcohol dependence.

LIMITATIONS

The sample size of 2400 might have led to a type II error, especially when
considering alcohol dependence.

Time to follow-up was short (18 months).

Cases of anxiety and depression that developed and resolved during
follow-up would not have been identified.

Acknowledgments

We thank the staff of the Office for National Statistics who were involved
in the fieldwork and data preparation of the Psychiatric Morbidity Among
Adults Living in Private Households, 2000 survey. The data collection was
funded by the Department of Health and the Scottish Executive Health
Department. However, the views expressed in this paper are those of the
authors alone and not necessarily those of the Department of Health or
Scottish Executive.